The E ect of Housing on Portfolio Choice Raj Chetty Harvard and NBER Adam Szeidl Central European University and CEPR October 2014 Abstract Economic theory predicts that home ownership should have a negative e ect on risktaking in nancial portfolios. However, empirical work has not found a strong relationship between housing and portfolios. We identify two reasons for the divergence between the theory and data. First, it is critical to distinguish between home equity wealth and mortgage debt, as they have opposite-signed e ects on portfolio choice. Second, it is important to isolate variation in home equity and mortgage debt that is orthogonal to unobserved determinants of portfolios. We estimate a model that permits home equity and mortgage debt to have di erent e ects on portfolio shares. We isolate plausibly exogenous variation in home equity and mortgages by using di erences across housing markets in average house prices and housing supply elasticities as instruments. Using data for 60,000 households, we nd that increases in property value (holding home equity constant) reduce stockholding signi cantly, while increases in home equity wealth (holding property value constant) raise stockholding. Our estimates imply that the stock share of liquid wealth would rise by 1 percentage point 6% of the mean stock share if a household were to spend 10% less on its house, holding xed wealth. We conclude that housing has substantial impacts on portfolio choice, as theory predicts. E-mail addresses: chetty@fas.harvard.edu and szeidla@ceu.hu. Thanks to Thomas Davido, Edward Glaeser, Albert Saiz, Stephen Shore, anonymous referees, and numerous seminar participants for helpful comments. Gregory Bruich, Calin Demian, Andras Komaromi, Jessica Laird, Keli Liu, James Mahon, Jeno Pal, Juan Carlos Suarez Serrato, and Philippe Wingender provided outstanding research assistance. We are grateful for funding from National Science Foundation Grants SES 0522073 and 0752835. Szeidl thanks the Alfred P. Sloan Foundation for support.
1 Introduction Houses are the largest assets owned by most households, but the impact of housing on nancial markets remains unclear. Theory predicts that housing generally reduces the demand for risky assets because it increases a household s exposure to risk and illiquidity (Grossman and Laroque 1990, Brueckner 1997, Flavin and Yamashita 2002, Chetty and Szeidl 2007). But empirical studies have not found a systematic relationship between housing and portfolios in practice (Fratantoni 1998, Heaton and Lucas 2000, Yamashita 2003, Cocco 2005). This paper reconciles the theory with the data. We identify two key factors, one theoretical and one empirical, that explain the discrepancy. Theoretically, we show that it is critical to separate the e ects of property value from the e ects of home equity to characterize the e ects of housing on portfolios. Empirically, we show that the endogeneity of housing choice biases prior estimates. Accounting for these two factors, we nd that increases in mortgage debt induce substantial reductions in the share of liquid wealth held in stocks, while increases in home equity wealth raise stock ownership. We structure our empirical analysis using a tractable model of portfolio choice that incorporates both the illiquidity and price risk e ects of housing. We rst characterize portfolio choice analytically in a stylized two-period model in which individuals move houses with an exogenous probability in the second period. We then show using numerical simulations that the key qualitative predictions of this model hold in richer environments that allows for endogenous moves, multiple periods, labor income risk, and stock market participation costs. The model predicts that property value and home equity have opposite-signed e ects on portfolio choice. Increases in property value (holding home equity wealth xed) generally reduce the stock share of liquid wealth by increasing illiquidity, increasing exposure to risk, and reducing the present value of lifetime wealth. In contrast, increases in home equity (holding property value xed) raise the stock share of liquid wealth with CRRA preferences through a wealth e ect, as emphasized by Yao and Zhang (2005). Since property value is the sum of mortgage debt and home equity, conditional on the level of home equity changes in mortgage debt are equivalent to changes in property value, and should reduce stockholding. However, a regression of portfolio shares on mortgage debt or on property value that does not fully control for wealth as in prior empirical studies may yield ambiguous estimates because the variation in property values could be driven by variation in home equity wealth. 1
Using the insights from the model, we turn to investigate the e ects of property value and home equity wealth on portfolios empirically. Because both portfolios and housing are endogenous choices that are a ected by unobserved factors such as background risk (Campbell and Cocco 2003, Cocco 2005), one cannot identify the causal e ect of housing on portfolios using cross-sectional variation across households. We use three research designs to address this central endogeneity problem. Each strategy isolates variation in mortgage debt and home equity that is orthogonal to unobserved determinants of portfolio choice under a di erent set of assumptions. Our rst research design instruments for property values and home equity using current and year-of-purchase home prices in the individual s state, calculated using repeat-sales indices. The current house price index is naturally a strong predictor of property values. However, the current house price also creates variation in a household s wealth: increases in house prices increase home equity wealth. To isolate the causal e ect of a more expensive house while holding wealth xed, we exploit the second instrument the average house price at the time of purchase. Individuals who purchase houses at a point when prices are high tend to have less home equity and a larger mortgage. We control for aggregate shocks and cross-sectional di erences across housing markets by including state and year xed e ects, thereby exploiting only di erential within-state variation for identi cation. We implement this cross-sectional IV strategy using microdata on housing and portfolios for 64,191 households from the Survey of Income and Program Participation (SIPP) panels spanning 1990 to 2004. We use two-stage-least-squares speci cations to estimate the e ect of property value and home equity on the share of liquid wealth that a household holds in stocks. We nd that housing has a large e ect on the share of stockholdings. A $10,000 increase in property value (holding xed home equity wealth) causes the stock share of liquid wealth to fall by 0.9 percentage points ($310), or 5.5% of mean stockholdings in the sample. This estimate is stable and statistically signi cant with p < 0:05 across a broad range of speci cations. In contrast, a $10,000 increase in home equity (holding xed total property value) increases the stock share of liquid wealth by 5.9% through a wealth e ect. 1 The elasticity of the stock share of liquid wealth with respect to outstanding mortgage debt is 1 To facilitate comparison between samples with di erent rates of stock market participation and hence di erent mean stock shares of liquid wealth, we report results in both percentages and percentage points throughout the paper. 2
-0.3, while the elasticity with respect to home equity wealth is 0.4. These portfolio changes are driven by both the extensive and intensive margins: changes in mortgage debt and home equity wealth induces changes in both the probability of owning any stocks and the amount of stocks held conditional on stock ownership. One potential concern with our rst research design is that state-level house price uctuations may be correlated with other factors such as local labor market conditions that directly impact portfolio choice. Our second research design addresses this concern. Here, we instrument for property values and home equity using the current and year-of-purchase national average of house prices interacted with the state housing supply elasticity, as measured by Saiz (2010) based on land availability and regulations. Intuitively, uctuations in the national housing market generate larger price uctuations in states with inelastic housing supply, generating di erential variation in house prices across states over time. This strategy yields estimates that are very similar to the rst design. We estimate that a $10,000 increase in property value causes a reduction in the stock share of liquid wealth of approximately 5.1%, while a $10,000 increase in home equity (holding xed total property value) raises the stock share by 4.7%. Our third research design uses panel data to study the short-run dynamics of portfolios from the year before to the year after home purchase. We test whether individuals who buy a larger house reduce their stock share of liquid wealth more than those who buy smaller houses. We again instrument for the change in property value using the state-level house price index at the time of home purchase. This panel strategy complements the cross-sectional approaches in two ways. First, it provides evidence that households actively change the composition of their nancial portfolios depending upon the amount they invest in a house. Second, it further mitigates concerns about the endogeneity of housing choices by permitting household xed e ects. Because the SIPP is a short panel, we observe portfolio shares both before and after home purchase for only 6,510 households. For this subset of households, we nd that a $10,000 increase in the price of the house leads to a 4.1% reduction in the stock share of liquid wealth in the year after home purchase, again very similar to the estimates from the rst two designs. This nding shows that stockholders primarily sell stocks (rather than bonds) to nance down payments. The magnitudes of the impact of housing on nancial portfolios can be assessed by consid- 3
ering various counterfactuals. First, suppose households have the same level of home equity wealth but spend 10% less on their house, so property value is 10% lower. The estimates from our the rst research design imply that the stock share of portfolios would be approximately 1 percentage point higher in this scenario. Given the mean level of liquid wealth in our sample of $40,000 (in 1990 dollars), this translates into a $400 increase in stockholdings per household on average. While this may appear to be a small change in absolute terms, it constitutes a 6% increase in the stock share of liquid wealth relative to the sample mean because many households do not hold any stocks. Among households that participate in the stock market, the predicted increase in the stock share from spending 10% less on housing is 4.6 percentage points. As an alternative counterfactual, suppose households have no mortgage debt and no home equity wealth. The net impact of having no housing wealth or liabilities would be an increase in the mean stock share of 5.2 percentage points (32%), or $2,100. Among stockholders, the share of liquid wealth held in stocks would increase by 19.9 percentage points. 2 Finally, as another metric, a one standard deviation increase in mortgage debt reduces the stock share of liquid wealth by 4 percentage points (25%). This is similar to the impact of a one standard deviation decrease in log nancial wealth on stock shares (Calvet, Campbell, and Sodini 2007). Our estimates of the e ect of housing on portfolios are larger and more robust than previous estimates. Fratantoni (1998) nds an elasticity of stock share with respect to mortgage debt of -0.15. In contrast, Heaton and Lucas (2000), Cocco (2005), and Yao and Zhang (2005) show that in cross-sectional OLS regressions in which property value is included as a covariate, the stock share is positively associated with mortgage debt. In related work, Yamashita (2003) nds an elasticity of stock share with respect to property value of approximately -0.1 (in a speci cation that does not include mortgage debt). Yamashita uses age, family size, home tenure, and aggregate housing returns as instruments for mortgage debt; unfortunately, these instruments are unlikely to be valid because standard models (e.g. Cocco 2005) generate direct relationships between all of these variables and portfolio choice, independent of the housing channel. Consistent with these prior studies, we also nd that OLS estimates in our data are often wrong-signed and are sensitive to covariates. Our IV 2 If the wealth taken out of housing were invested in other assets (e.g., nancial assets) so that total wealth remained xed, the stock share of liquid wealth would likely rise even further. 4
estimates are less sensitive to speci cation because they are driven by variation that is orthogonal to most household-level determinants of portfolios and because we systematically separate the e ects of mortgage debt and home equity. The robustness of these results is underscored by a recent study by Fougere and Poulhes (2014), who replicate our analysis using data on French households and con rm that when one isolates exogenous variation in these variables, mortgage debt and home equity have signi cant, opposite-signed e ects on portfolio shares. The link between housing and nancial decisions that we document here has implications for several issues. For example, our results suggest that recent increases in leverage due to the easing of credit in the U.S. (Mian and Su 2011) may have induced households to withdraw funds from the stock market. This reduction in demand for risky assets could have further precipitated the sharp decline in asset prices. Our results also suggest that homeownership ampli es the welfare cost of shocks. Policies such as unemployment and health insurance or restrictions on the riskiness of nancial portfolios could therefore generate signi cant welfare gains for individuals who own houses or other risky, illiquid assets. The remainder of the paper is organized as follows. The next section presents a portfolio choice model, analyzes its comparative statics with respect to housing, and quanti es the impacts one should expect using numerical simulations. Section 3 describes the data. Section 4 presents the empirical results. Section 5 concludes. 2 Theoretical Predictions In this section, we characterize the forces through which exogenous changes in property value and home equity a ect household portfolios. We begin by deriving an approximate analytical expression for optimal portfolio shares in a stylized two-period model. This stylized model provides a simple, tractable framework that uni es the intuitions of several papers that have highlighted di erent mechanisms through which housing a ects portfolio choice, including illiquidity (Grossman and Laroque 1990, Chetty and Szeidl 2007), home price risk (Flavin and Yamashita 2002), hedging e ects (Sinai and Souleles 2005), and diversi cation e ects (Yao and Zhang 2005). We then generalize the model to allow for xed moving costs, stock market participation costs, labor income risk, and dynamics. Using numerical simulations, we show that the key comparative statics of the two-period model hold with plausible parametrizations 5
in these extensions. 2.1 Stylized Two-Period Model Our stylized model builds on Cocco s (2005) model of housing and portfolio choice, which incorporates all of the mechanisms described above but does not permit an analytical solution. To obtain an analytic expression for portfolio shares, we make a number of simplifying assumptions, most importantly that households can only move at exogenous random dates. The more realistic model in which households can move by paying a xed cost is analytically intractable, and we therefore analyze it using numerical methods below. A household endowed with a house H 0, mortgage debt M 0, and liquid wealth L 0 makes a nancial portfolio investment decision in t = 0. Consumption takes place in t = 1, and the household maximizes E 0 h i 1 C 1 1 H 1 1 (1) where C 1 is adjustable (e.g., food) consumption and H 1 is housing consumption. As in Campbell and Cocco (2003), we assume that moves in t = 1 are exogenous. With probability the household stays in the current house (H 1 = H 0 ), while with probability 1 it moves, and chooses H 1 optimally. One interpretation of this assumption is that the xed cost of moving is su ciently high that except for life-changing events, such as marriage or childbirth, the household does not consider changing houses when making nancial investments. In this model, measures the strength of housing commitment. At t = 0 the household can invest in a riskfree nancial asset with return 1+R f = exp (r f ) and a risky asset with return 1 + R = exp (r), where r is normally distributed with mean r and variance 2 r. The only choice variable at t = 0 is, the share of the risky asset out of liquid wealth. Let R p = R + (1 ) R f denote the household s nancial return, and assume that short sales constraints restrict 2 [0; 1]. Home prices are P 0 = 1 and P 1 = exp (p 1 ), where p 1 is normal with mean p and variance 2 p. The correlation between home price growth and stock returns is = corr[p 1 ; r]. The household chooses to maximize (1) subject to the budget constraint C 1 + P 1 H 1 = (1 + R p ) L 0 + Y 1 + P 1 H 0 (1 + R m ) M 0 6
where R m is the mortgage rate and Y 1 is labor income, which for now we assume is deterministic. Let the (risk-adjusted) present values of mortgage debt, labor income, liquid wealth, home value, and lifetime wealth be denoted by M = M 0 (1 + R m ) = (1 + R f ), Y = Y 1 = (1 + R f ), L = L 0, P H = P 0 H 0, and W = L + Y + P H M. Optimal portfolio shares. share using log-linearization. equation: r We derive an approximate equation for the optimal stock Household optimization yields the following log-linear Euler r f + 2 r 2 = cov r; vnm 0 + (1 ) cov r; vm 0, (2) where v 0 nm and v 0 m are the log marginal utilities of wealth in t = 1 in the no move and move states of the world and the weight = 1 + 1 1 : (3) (1 ) (1 ) + (P H=W ) (1 ) (1 P H=W ) + The intuition for (2) is that the agent optimizes by trading o the expected gain from investing in the risky asset with the additional uctuation in marginal utilities he bears as a result of the investment. The additional risk is measured by the covariance of the market return with marginal utilities, weighted by. The weight can be interpreted as a marginal-utilityadjusted probability of not moving, analogous to a state-price density. When the housing share of lifetime wealth P H=W equals the optimal share, equation (3) implies that =. But when P H=W >, > : since the household starts with too much housing and too little adjustable consumption, the marginal utility of wealth is on average relatively higher in the no-move state. As a result, the consumer is more sensitive to uctuations in this state, explaining the larger weight. An approximation for the optimal portfolio share can be derived from the Euler equation using standard methods (see e.g., Campbell and Viceira 2002): Proposition 1 Letting c = + t = 0 is, to a log-linear approximation,, the optimal share of stocks out of liquid wealth at = 2 r r r f + h 2 r=2 i + cov [p 1 ; r] (1 ) c L W P H + (1 ) L W 2 r ( 1) P H h W i: c L W P H + (1 ) L W (4) 7
Proof: See Appendix. To understand this expression, rst consider the case in which house prices do not covary with stock prices (cov[p 1 ; r] = 0). In this case, the second term drops out and (4) has an interpretation analogous to a familiar myopic rule: the numerator measures the expected excess return of stocks, while the denominator equals stock market risk 2 r multiplied by e ective risk aversion over liquid wealth. Because housing is a xed commitment, risk aversion is the weighted average c L= (W P H)+(1 ) (L=W ). When the consumer is free to move ( = = 0), this term simpli es to L=W, yielding the classic Merton (1969) formula adjusted for the fact that stocks are measured as a share of liquid rather than total wealth. When the consumer can never adjust housing ( = = 1), e ective risk aversion is c L= (W P H). This e ective risk aversion is di erent for two reasons. First, because the agent cannot move, shocks are concentrated on adjustable consumption W P H and hence have an ampli ed e ect on marginal utility (Chetty and Szeidl 2007). Second, because H 1 does not adjust, curvature is determined by (1 ) (1 ) in (1), generating the c term. Finally, when cov[p 1 ; r] 6= 0, home price risk generates a hedging demand for stocks, re ected in the second term in (4). This term is also a ected by the strength of the housing commitment through. When = = 1, the home is never sold, and hence home price risk does not a ect behavior (Sinai and Souleles 2005). Comparative Statics. We are interested in characterizing how the optimal portfolio share varies with property value P H and total wealth W. With CRRA utility, the household seeks to maintain a constant share of its total wealth in risky assets as W rises. Therefore, an exogenous increase in home equity wealth which is relatively safe induces the household to buy more stocks. This diversi cation e ect (Yao and Zhang 2005) is captured by the terms involving W in the denominator in (4). In our model, an increase in wealth also reduces, the weight of the no-move state in the Euler equation. Because the no-move state is typically riskier, this additional e ect generally acts to further raise. Exogenous increases in property value P H reduce through three channels. First, for a given W, increasing P H implies that a larger share of wealth is tied up in housing, making marginal utility higher and more sensitive to shocks in the no-move state. This e ect arises from an increase in e ective risk aversion c L= (W P H) in the denominator of (4) and by a higher weight on the no-move state. Second, when cov[p 1 ; r] > 0, a higher P H results in 8
greater exposure to home price risk, which has a negative e ect on hedging demand. Third, holding xed home equity, a higher property value means higher mortgage debt. If the mortgage rate exceeds the risk free rate (R m > R f ), increased mortgage payments reduce lifetime wealth W, resulting in lower stockholdings in (4). These comparative statics show that it is critical to distinguish changes in property value from changes in home equity wealth to uncover the e ects of housing on portfolio choice. Increases in property value that come from more mortgage debt reduce stockholding, while increases in property value that are accompanied by additional home equity wealth have ambiguous e ects. 2.2 Numerical Results in the Static Model We now assess the quantitative importance of the comparative statics and evaluate their robustness to incorporating additional features into the stylized model. In this subsection, we extend the model to incorporate (1) xed adjustment costs, which permit households to move at any time by paying a cost, (2) stock market participation costs, and (3) labor income risk. Because these features make the model analytically intractable, we use numerical methods to characterize the relationship between housing and portfolios. 3 We begin by calibrating the parameters of the model based on the existing literature. For parameters related to life-cycle portfolio choice, we follow Cocco, Gomes, and Maenhout (2005) and set the annual riskfree return at R f = 0:02, the annual equity premium at ER R f = 0:04, the standard deviation of the log stock return at = 0:157, and the coe cient of relative risk aversion at = 10. For housing related parameters, we set the relative preference for housing at = 0:2 (Yao and Zhang 2005), the annual mean growth and standard deviation of home prices at p = 0:016 and p = 0:062, and the annual mortgage rate at R m = 0:04 (Cocco 2005). Both Cocco and Yao and Zhang assume a zero correlation between housing and the stock market. To get a sense of the e ect of this correlation which a ects the riskiness of housing on portfolios, we report results with both = 0 and = 0:1. We set the time horizon of our static model to balance two forces. First, because the average age in the sample we study below is 48 years, households have approximately 30 3 Details on the solution methods are given in the appendix. We thank Joao Cocco for sharing his code for solving the model in Cocco, Gomes, and Maenhout (2005). 9
years as their investment horizon. Second, because the household is likely to repay its mortgage and move with high probability over the course of multiple decades, the relevant horizon for studying housing commitments is signi cantly shorter. Because our static model cannot simultaneously handle both of these horizons, and given that our main interest is the e ects of housing commitments, we set the length of time between t = 0 and t = 1 to be 10 years. Importantly, we use this horizon simply as a benchmark; solving the model with longer horizons yield qualitatively similar results. In addition, in the next subsection we study a dynamic model with a twenty year horizon that more fully incorporates both the short-term and the long-term factors that a ect portfolio choice. Cocco (2005) estimates a moving probability of 24.4% over a ve year horizon, which implies that the probability a household does not move over a ten year horizon is 57%. We therefore set a baseline = 0:55, but evaluate the sensitivity of our results to other values as well. We assume that the household has liquid wealth L 0 = $40; 000, home value P 0 H 0 = $125; 000 and mortgage M 0 = $53; 000, the sample means in our data (see Table 3a below). Cocco, Gomes, and Maenhout (2005, Figure 3b) report that the ratio of the riskadjusted present value of future labor income to current nancial wealth is approximately 5 for households in their late forties and early fties. We therefore set labor income at Y 1 = 5L 0. Numerical solution of stylized model. We rst report the numerical solution of the stylized two-period model as a reference to verify that the approximate solution in (4) accurately captures the comparative statics of the model. 4 Table 1 reports optimal portfolio shares as a function of property value and home equity for a range of model parameters. Panel A con rms that increases in property value P 0 H 0 (holding xed home equity wealth) reduce the optimal share of stocks. For example, when the probability that the household does not move is = 0:55 and the correlation between housing and returns is = 0, increasing property value from $125; 000 to $135; 000 results in a reduction in stock share from 66:5% to 60:1%, or by about 9:6%. 5 Panel B considers the e ect of changes in home equity wealth 4 The quality of the approximation is high for short horizons but deteriorates slightly over longer horizons. For example, the average absolute di erence between the numerical and the approximate solution across all the parameter values considered in Table 1 for a one-year horizon is only 0.05 percentage points. With a ve year horizon, the mean error grows to 0.34 percentage points and for ten years it is 2.18 percentage points. Despite these deviations, the approximate solution shows the same patterns as the numerical results. 5 Our calibrated model produces stock shares that are substantially higher than the mean stock share 10
(holding xed property value). For the same parameters, an increase in home equity from $72; 000 to $82; 000, while holding home value xed at $125; 000, increases the stock share from 66:5% to 72:6%, or by about 9:2%. We observe qualitatively similar e ects for other parameter values. Having established the comparative statics of interest in our stylized model, we now consider a series of generalizations. Fixed adjustment costs. We begin by relaxing the assumption that households can only move at random, exogenous dates. A more realistic assumption is that households can move at any time by paying a xed cost. Let denote the size of this xed cost as a share of property value. Smith, Rosen, and Fallis (1988) estimate the monetary component of moving costs to be approximately = 0:1. We consider values of = 0:1 and = 0:2, the latter of which captures other utility costs of moves (e.g., the need to change a child s school). Panel A of Table 2a reports results analogous to those in Table 1 from this model. The direction of comparative statics are generally the same, although the property value e ects are smaller in magnitude, as should be expected given that housing is a weaker commitment in this model. One interesting feature of the xed cost model is that the comparative statics of interest change sign for some parameter values. For instance, when = 0:2 and = 0, increasing home value from $105; 000 to $115; 000 increases the stock share from 84:2% to 85:2%. Additional increases in home value beyond this level reduce the stock share as in the exogenous moves model. Such non-monotonicities in risk preferences in the presence of xed costs were rst observed by Grossman and Laroque (1990) and more extensively documented by Yao and Zhang (2005). The intuition is that households who are relatively close to the boundary of their inaction region have a gambling motive: by holding more stocks, they can increase the probability of buying their ideal house. For households who are on the margin of moving, this mechanism can sometimes overpower the three forces that act toward reducing. Table 2a shows that for most parameter values, the other three forces dominate. However, the fact that the model can sometimes produce a positive relationship implies that the average e ect of property value on the stock share is ultimately an empirical question. of liquid wealth in our data (16%). The model matches the data better if we focus on the subsample of stockholders, among whom the average stock share is 55%. As Cocco, Gomes and Maenhout (2005) and Davis, Kubler and Willen (2006) emphasize, calibrated models of portfolio choice frequently overpredict stockholdings because relatively safe future labor income creates an incentive to leverage nancial investment. As we discuss below, xed costs of stock market participation can help address this issue. 11
Participation costs. Next, we consider the e ect of incorporating stock-market participation costs. We extend our baseline model by assuming that the household must pay a xed cost F at t = 0 if it wishes to hold any stocks. Vissing-Jorgensen (2003) estimates that, when allowing for cross-sectional variation in xed costs, a cost distribution with median cost of $350 per year (in 1982-84 prices) can explain the pattern of non-participation in 1994. Converting this median estimate to 1990 prices (the units we use to measure wealth and home value in our empirical analysis) and computing the present value of paying this amount every year for ten years to re ect the investment horizon, we obtain an estimate of F = $4; 207. Panel C of Table 2a reports optimal portfolios in the presence of this cost with = 0:55 and = 0. Our model predicts an active extensive margin: for example, as property value increases from $135; 000 to $145; 000, the household changes the stock share from 64:8% to zero. Intuitively, a higher property value leads the household to reduce the stock share, but the xed cost of participation outweighs the bene t of investing a small amount in stocks, inducing the household to exit the stock market entirely. These extensive margin responses amplify the e ects of housing on portfolio shares, but the qualitative predictions of the model remain similar. 6 Income risk. Next, we consider the e ects of labor income risk by allowing labor income Y 1 to be stochastic. Because the household must repay the exogenously xed mortgage (i.e., there is no default), labor income must be bounded from below for the model to be well-de ned. We thus assume that Y 1 = Y1 s + Y 1 r where both terms are non-negative. Y s 1 is a safe (deterministic) component of labor income, while Y r 1 In keeping with the earlier parametrization, we assume EY 1 is a lognormal random variable. = Y s 1 + EY r 1 = 5L 0. We set (somewhat arbitrarily) the safe share of expected labor income to be 60%: Y s 1 = 3L 0 and EY r 1 = 2L 0. We assume that Y r 1 and P 1 are jointly lognormal, and set V ar [log (Y r 1 )] and Cov [log (Y r 1 ) ; log (P 1)] to match the standard deviation of log (Y 1 ) and the correlation between log (Y 1 ) and log (P 1 ). We calibrate log (Y 1 ) to the annual standard deviation of 0.13 for labor income growth as used by Yao and Zhang (2005). For the correlation between log (Y 1 ) and log (P 1 ), we consider both zero (as a benchmark) and 0.55 (as in Cocco, 2005). 6 Vissing-Jorgensen (2003) also reports a lower xed cost estimate of $150 per year (in 1982-84 prices), which explains three quarters of non-participation in a framework that does not permit cross-sectional variation in the xed cost. For the parameters considered in Table 2a, this lower xed cost does not generate extensive margin responses, as the household always chooses to participate in the stock market. 12
As above, we assume = 0:55 and = 0. Panel C of Table 2a shows that introducing labor income risk reduces the stock share. Intuitively, these shocks increase background risk and hence reduce the risk appetite of investors. However, the predictions about the e ect of property value and home equity are again unchanged in sign and remain similar in magnitude. The correlation between labor income and home prices has small e ects on portfolio shares. This is likely because home price risk () itself has small e ects. Intuitively, house prices only matter in the event of a move, and even then, much of the money invested in the previous house is used to purchase the new house, providing a natural hedge against house price risk (Sinai and Souleles 2005). 2.3 Dynamic Extensions The model we have analyzed thus far is e ectively static: there is a single decision about portfolio choice, and all uncertainty is resolved in a single period. We now consider three extensions that make the model dynamic: (1) consumption in the initial period, (2) persistent uncertainty, and (3) a bequest motive. To isolate the e ect of these changes, we consider each separately. Consumption in the initial period. To allow for consumption and savings decisions in the initial period, suppose that at t = 0 the household can freely choose adjustable consumption C 0. Housing consumption is restricted to be equal to the housing endowment H 0. Preferences are now given by h i 1 C 1 0 H 0 1 + E 0 h i 1 C 1 1 H 1 1 : We also assume that the household receives initial income Y 0 in period zero. We set = 0:55 and = 0, use an annual discount factor of 0:96, and set Y 0 such that the household with property value $125; 000 and home equity $72; 000 saves exactly the same amount ($40,000) as the exogenous starting level of liquid wealth we used in the static model above. This gives Y 0 = $190; 195, which is slightly lower than Y 1. The qualitative e ects of property value and home equity in this setting are the same as in the static case. However, the stock share is now more sensitive to changes in property value and home equity. This is because of the additional margin of savings. For example, 13
when higher property value increases future risk, the household not only reduces the dollars it invests in stocks (as in the baseline model), but also saves more, further lowering the share of stocks in liquid wealth. Persistent uncertainty. To consider the e ects of multi-period uncertainty, we now introduce a third period into the baseline model. In t = 0, the household makes a portfolio decision as in the baseline version of the model. In t = 1, it repays its outstanding mortgage, earns labor income Y 1, moves houses with (exogenous) probability (1 ), consumes, and makes a new portfolio decision. In t = 2, the household earns labor income Y 2, moves with independent probability (1 ), and consumes. We assume that each period lasts ten years. We set Y 1 = 4L 0, Y 2 = 3L 0, = 0:55, and = 0. The annual discount factor is 0:96. Panel B of Table 2b shows that in this environment with persistent uncertainty, the e ects of property value and home equity, though larger in magnitude, are once again qualitatively similar to the baseline speci cation. Thus, the results of the one-period model with a ten year horizon continue to serve as a useful benchmark. Bequests. Finally, we address the concern that the household cannot monetize the house at the end of t = 1 in our baseline model by introducing a bequest motive. Following Cocco (2005), we assume that the household bequeaths the house as well as any unconsumed savings (S 1 ) to its o spring, who derive CRRA utility from the total market value of these assets. Thus implicitly we assume that death is a move-inducing event. Similarly to Cocco (2005), total utility from the perspective of t = 0 is given by E 0 h i 1 C 1 1 H 1 1 [P 1 H 1 + S 1 ] 1 + E 0 : 1 Panel C of Table 2b reports the results from this speci cation. The qualitative results remain similar, but now the stock share responds less to changes in property value and home equity. One force that explains this pattern is that bequest utility is bounded from below due to the presence of housing. Intuitively, because parents know that their children will have the house even if they cannot bequeath any nancial assets, they are less sensitive to changes in risk. Together, the results in Tables 1 and 2 show that mortgage debt generally reduces stockholding while home equity wealth increases it. However, the magnitudes of these e ects are sensitive to model speci cation. The quantitative impacts of housing on portfolio choice are 14
therefore an empirical question, to which we now turn. 3 Data and Sample De nition We estimate equation the e ects of housing on portfolio choice using data from seven Survey of Income and Program Participation panels that began in years 1990-2001. Each SIPP panel tracks 20,000 to 30,000 households over a period of 2-3 years, collecting information on income, assets, and demographics. During the rst four panels, asset data were only collected once; in the last three panels, asset data were collected once per year, permitting a panel analysis of changes in portfolios. The main advantages of the SIPP relative to other commonly used datasets on nancial characteristics such as the Survey of Consumer Finances are its large sample size and detailed information about covariates such a complete housing history and geographic location. We obtain quarterly data on average of housing prices by state from 1975-2004 using the repeat sales index constructed by the O ce of Federal Housing Enterprise Oversight (OFHEO). Calhoun (1996) provides a detailed description of the construction of the OFHEO index, which has been widely used in studies of housing markets (see e.g., Himmelberg, Mayer, and Sinai 2005). 7 We obtain land topology-based measures of housing supply elasticity by state from Saiz (2010). 8 Saiz predicts housing supply elasticities using data on physical and regulatory constraints (land availability and use regulations), providing a convenient index of the supply constraints in each housing market. The seven SIPP panels together contain information on 163,405 unique households, of which 97,798 are homeowners, whom we de ne as individuals with positive property value and positive home equity. 9 70,924 of these households bought their current house after 1975 7 We use OFHEO price indices rather than other popular measures such as Case-Shiller indices because Case-Shiller data are available only starting in 2000 for selected metro areas. Unfortunately, geographic information below the state level is not available for more than two-thirds of the observations in our sample. Although the two indices di er in the way they treat appraisals and the set of loans they consider, Leventis (2007) reports a correlation of 0.98 between the OFHEO and Case-Shiller indices for markets where both measures are available. 8 We aggregate the MSA-level statistics reported by Saiz (2010, Table 6) to the state level by taking population-weighted means across MSAs within each state. For MSAs that cross state boundaries, we use the population in the MSA within each state, calculated from Census statistics on county population. 9 4.36% of households with positive property value report zero or negative home equity. We exclude these individuals in our baseline analysis because we use log home equity as an independent variable in some speci cations and wish to retain a xed sample across all speci cations. Including these households has negligible impact on the estimates in levels; for instance, for the speci cation in Column 4 of Table 5, we nd 15
and therefore have OFHEO data for the year of home purchase, which is required for our instrumental variable analysis. We exclude an additional 6,733 households whose reported liquid wealth by our de nition is zero, making their portfolio shares ill de ned. These exclusions leave us with 64,191 homeowners in our cross-sectional analysis sample. Table 3a reports summary statistics for the cross-sectional analysis sample. 10 In the cross-sectional sample, homeowners own houses that are worth approximately $125,000 on average in 1990 dollars. The average amount of home equity is $72,000 and the average outstanding mortgage is $53,000. The average household head is 48 years old and has lived in his current house for 8.4 years. Mean total wealth (which includes liquid wealth, home equity, wealth in retirement accounts, and other illiquid assets such as cars) is $173,000. 11 We de ne liquid wealth as the sum of assets held in stocks, bonds, checking, and savings accounts, excluding retirement accounts. 12 Mean liquid wealth is $40,000, but this distribution is very skewed; the median level of liquid wealth is only $5,600. 13 Households hold on average approximately 16% of their liquid wealth in the form of stocks in taxable (non-retirement) accounts and 84% in safe assets (bonds, checking, and savings accounts). The relatively small fraction of wealth held in stocks re ects the fact that only 29% of the households in the data hold stocks outside their retirement accounts, consistent with Vissing-Jorgensen (2003). Panel data on portfolio shares are available for households in the 1996 and 2001 SIPP panels. In these panels, data on portfolio shares were collected annually, giving us information on assets and homeownership between 3 to 4 times per household. We form our panel analysis sample using the 6,150 observations for which we observe a purchase of a new house within the panel and have data on portfolio shares both before and after this home purchase. 14 a coe cient of -8.19% (s.e. 2.97%) on property value and 8.95% (s.e. 3.33%) on home equity if we include these households. 10 See Appendix Table 1 for summary statistics for the full SIPP sample. 11 Total wealth is gross household wealth measured on the survey. It includes nancial assets as well as all real estate (including second homes), cars, and private business equity. Debts are not subtracted from the total wealth measure. 12 We exclude retirement accounts from our de nition of liquid wealth because households typically incur signi cant penalties to withdraw money from retirement accounts prior to retirement. Moreover, the SIPP does not contain data on portfolio allocations within retirement accounts, so we cannot study changes in portfolio choice behavior within these accounts. 13 Skewness and outliers do not a ect the results reported below. Trimming outliers (e.g. by excluding the top and bottom 1% of households by wealth or property value) has virtually no e ect on our 2SLS estimates. This is because the distribution of predicted housing values generated by the instruments is not skewed. There are few outliers in the tted values from the rst stage. 14 When we include these households in the cross-sectional sample, we only use data from the rst year in which assets are observed. Hence, each observation in the cross-sectional sample is for a unique household. 16
Table 3b reports summary statistics for the sample we use in the panel analysis. Homeowners in the panel sample generally have similar characteristics to those in the cross-sectional sample, with three exceptions. First, they have less home equity and more mortgage debt, as expected for new home buyers. Second, they are slightly less wealthy, consistent with being younger on average. Finally, they hold more stocks in their portfolios. This is because the panel sample spans 1996-2003, a period with higher stock ownership than the early 1990s. 4 Empirical Analysis We estimate the impacts of property value and home equity using the following linear speci- cation for portfolio shares: stock share i = const + 1 property value i + 2 home equity i + X i + " i (5) where X i denotes a vector of controls, including components of total wealth such as liquid wealth and income. The model in Section 2 predicts 1 < 0 and 2 > 0. 15 The error term " captures other sources of heterogeneity in portfolios. These may include entrepreneurial risk (Heaton and Lucas 2000), investment mistakes (Odean 1999, Calvet, Campbell and Sodini 2007), heterogeneity in risk aversion, or measurement error in income (Cocco 2005). Some of the e ects captured by the error term may be correlated with property value, creating bias in OLS estimates of 1 and 2. For instance, Cocco (2005) emphasizes biases due to unobserved labor income, which a ects both the stock share and property value. Suppose that Y 1 = Y obs 1 +Y un 1 where only Y obs 1 is observed to the econometrician. Since higher lifetime wealth generates higher stockholdings, " is positively related to Y un. If households with higher future labor income own larger houses as predicted by the model with persistent uncertainty property value is also positively related to Y un, and hence the OLS estimate of 1 is biased upward. Indeed, Cocco (2005, Table 6) runs cross-sectional OLS regressions using simulated data from his model and nds a positive e ect of mortgage debt on stockholdings 15 An alternative speci cation is to normalize the housing variables by liquid wealth. We show that our results are robust to such a speci cation, but opt to use levels in our baseline model for two reasons. First, when liquid wealth is imperfectly measured and close to zero for some observations, normalizing by it introduces large outliers in the independent variables of interest. Second, our simulations show that one should nd a relationship between the stock share and levels of property value and home equity wealth. 17
caused by omitting future labor income from the regression. Such endogeneity problems make it essential to isolate variation in property value and home equity that is orthogonal to " in order to identify 1 and 2. We divide our empirical analysis into four sections. First, we con rm that estimating (5) using OLS in our data yields results that are similar to those of prior studies. We then identify the causal impacts of mortgage debt and home equity wealth on portfolios by using three di erent research designs to estimate (5): variation in mean house prices, variation in local housing supply constraints, and changes in portfolio shares around home purchase in panel data. 4.1 OLS Estimates Previous studies have estimated OLS regressions of portfolio shares on property values, mortgage debt, and home equity with various control vectors and obtained mixed results. To ensure that the di erences between our ndings and theirs are not driven by di erences in data or sample de nitions, we begin by estimating similar speci cations in our sample. Column 1 of Table 4 reports OLS estimates of a regression of the stock share of liquid wealth on property value and home equity wealth without any covariates. Consistent with the ndings of Heaton and Lucas (2000), Cocco (2005), and Yao and Zhang (2005), we nd that an increase in property value (mortgage debt) is positively associated with the stock share of liquid wealth, contrary to the model s predictions. This is presumably because individuals with larger properties tend to be wealthier or face less background risk and these omitted factors induce them to hold more stocks. In column 2, we attempt to account for some of these factors by including controls for household income and private business wealth; household head s education, number of children, and age; and a 10 piece linear spline in liquid wealth to control exibility for a household s level of wealth. The inclusion of these covariates reduces the coe cient on property value by approximately 80%, but it remains positive in sign. In column 3, we exclude households with zero mortgage debt, who constitute 23% of homeowners in the sample, as in Fratantoni (1998). This change in sample speci cation makes the coe cient on property value negative and statistically signi cant, consistent with Fratantoni s ndings. Importantly, Fratantoni is not able to control for location as the SCF 18
does not contain geographic information. Once indicators for state of residence are included, the negative correlation between property values and stock shares is no longer signi cant, as shown in Column 4 of Table 4. These OLS results echo the instability of estimates found in prior studies. Moreover, they indicate that the endogeneity of housing choices is likely to bias the e ect of property value on stock shares upward. These ndings call for research designs that isolate variation in mortgage debt and home equity that is less correlated with unobserved determinants of portfolios. 4.2 Research Design 1: Mean House Prices Identi cation Strategy. Our rst research design exploits two instruments to generate variation in home equity and property value: the average price of houses in the individual s state in the current year (the year in which portfolios are measured) and the average price of houses in the individual s state in the year that he bought his house. The intuition for this identi- cation strategy is illustrated in Figure 1, which plots average real home prices in California from 1975-2005 using the OFHEO data. Consider a hypothetical experiment involving a set of individuals who buy identical houses and only pay the interest on their mortgage (so that debt outstanding does not change over time). As a baseline, consider individual A who buys a house in 1985 (dashed red line) and whose portfolio we observe in 2000 (solid blue line), as shown in Panel A. Now compare this individual to individual B who buys the same house in 1990 and whose portfolio we also observe in 2000. Individuals A and B have the same current property value, but individual B is likely to have less home equity and a larger mortgage, because home prices were higher in 1990 than 1985. Intuitively, since individual B is buying the same house at a higher price, he needs a bigger mortgage; and because he enjoys less home price appreciation than A, he will end up with lower home equity in 2000. Now consider a second experiment, comparing panel C to A. Individual C buys the same house in 1985, but we observe his portfolio in 2005. This individual has the same mortgage debt as individual A (under the assumption that individuals only pay interest to service their debt), but has higher home equity and wealth at the time we observe his portfolio. Together, the two experiments (instruments) allow us to separately identify the causal e ects of mortgages and 19